The Family Stress Model (FSM: Conger et al., 1992) was first devised to understand better the impact of negative economic events on families in the Midwestern United States during the Great Farm Crisis in the 1980s. The earliest FSM studies used samples of White families in rural farming communities in Iowa (Conger et al., 1990, 1992, 1993, 1994) and showed that negative economic events were associated with poor outcomes for children mainly through their effects on parents’ mental health, relationship quality, and parenting behaviors. Specifically, the FSM posits that economic pressures arising from negative economic events such as low family income, income loss, unstable work, or debts can lead to higher levels of depressive moods for both mothers and fathers, which then lead to relationship strain in the form of interparental conflict. Subsequently, poor interparental relationship quality is linked to lower involved or nurturant parenting behaviors that ultimately result in children’s maladjustment (Conger et al., 1992).

Expanding on this work, researchers have also tested the FSM with racially diverse samples and have found support for the model (Conger et al., 2002; Parke et al., 2004; Masarik and Conger, 2017; Gard et al., 2020; Curran et al., 2021; Lee et al., 2021). For example, Lee et al. (2021) recently applied the FSM to a sample of BSF families and found that fathers’ depressive symptoms was a mediating path between material hardship, but not income poverty, and destructive interparental conflict. Curran et al. (2021) also applied the FSM to a BSF sample and showed in cross-lagged panel models that fathers’ depressive symptoms at the 15-month follow-up predicted higher levels of destructive interparental conflict at the 36-month follow-up, but not vice versa. Both studies underscore the centrality of paternal mental health as a significant factor affecting family processes, namely interparental relationship quality.

Neither BSF study, however, included parenting nor child outcomes, and more importantly, both focused on testing the FSM looking at family conflict and poor mental health and did not use a risk and resilience framework and consider the buffering effects of positive family dynamics. The current study was designed to test how a supportive coparenting alliance between mothers and fathers predicted responsive parenting and in turn, children’s prosocial behavior in an effort to look at protective factors within families experiencing material hardship.

Prior studies have examined the links between material hardship, coparenting alliance, and responsive parenting behaviors (Gershoff et al., 2007; LeBaron et al., 2020; Curran et al., 2021). Coparenting alliance is often characterized by both parents’ investments in their children, a respect for each other’s judgment about child rearing, and a desire to communicate child-related information (Weissman and Cohen, 1985; Feinberg, 2003). Recently, LeBaron et al. (2020) used a sample of two-parent families from the BSF project and showed that material hardship at the 15-month follow-up was linked with lower levels of fathers’ perceived coparenting alliance (i.e., communication, support, and teamwork), but not mothers’ perceived coparenting alliance, at the 36-month follow-up. The researchers noted the possibility that when fathers with low income are faced with financial strain that makes it difficult to help meet their families’ material needs, they may end up prioritizing financially providing for their families over building a coparenting alliance with mothers (LeBaron et al., 2020). That is, stress with meeting their families’ material needs may undermine socioeconomically disadvantaged fathers’ abilities to successfully engage in positive coparenting behaviors with their partners. Alternatively, mothers may be more likely to engage in gatekeeping behaviors when fathers do not meet breadwinner norms (e.g., unemployed) (Waller, 2012) and the financial stress associated with material hardship and meeting the needs of the family may take its toll on the coparenting relationship. Unlike LeBaron et al. (2020) though, Curran et al. (2021) in their cross-lagged modeling of material hardship and coparenting alliance using BSF data found that material hardship at the 15-month follow-up was not associated with either mothers’ or fathers’ perceived coparenting alliance at the 36-month follow-up.

Findings on material hardship and responsive parenting also seem to be mixed, and available studies seem to primarily focus on mothers. In one study examining links between material hardship and mothers’ positive parenting, Shelleby (2018) used data from the Fragile Families Child Wellbeing Study (FFCWS) and found that material hardship when children were a year old was not linked with mothers’ positive parenting (e.g., praise child, warmth) when children were 5 years old. They did not include information on fathers, even though work cited earlier suggested that men’s mental health was a contributing factor to family conflict. Gershoff et al. (2007) also focused predominantly on mothers by using data from the Early Childhood Longitudinal Study-Kindergarten cohort (ECLS-K) and found that material hardship was linked with higher levels of maternal positive parenting (e.g., warmth, cognitive stimulation) when children were 6 years old—a finding that was unexpected. The researchers noted that mothers may be investing in positive parenting behaviors, when they are unable to provide economic resources to improve their children’s lives. Few studies focus specifically on material hardship and fathers’ positive parenting, and instead use indicators of fathers’ economic conditions (e.g., employment status, living in poverty) to examine relations between fathers’ parenting and children’s outcomes (Johnson, 2001; Waller, 2012; Baker et al., 2018). For example, using a sample of fathers from the FFCWS, Waller (2012) showed that fathers being employed when their children were 3 years old was associated with mothers’ reports of fathers spending more time with their children but fathers engaging in a lower number of daily activities (e.g., playing outside, reading stories, and singing songs).

When studies do include both mothers and fathers from socioeconomically disadvantaged backgrounds, which again are limited in number, there is evidence that lack of economic resources can negatively affect the quality of parent-child relationships. For instance, Baker et al. (2018) examined fathers and mothers from the Early Childhood Longitudinal Study-Birth Cohort (ECLS-B) and showed that poverty levels were related to lower levels of paternal warmth and cognitive stimulation during fathers’ interactions with their 24-month-old children in the home. Family poverty was associated only with lower levels of cognitive stimulation during mother-child interactions. Overall, given the mixed results of prior research and limited number of studies including both mothers and fathers, additional research is needed to understand better the links between material hardship, the coparenting alliance, and mothers’ and fathers’ responsive parenting among families with low income.

Research has examined relations specifically between the coparenting alliance and positive parenting behaviors for both mothers and fathers from socioeconomically disadvantaged backgrounds (Jones et al., 2005; Shook et al., 2010; Barnett et al., 2011; Fagan and Palkovitz, 2011; Lee et al., 2020). For example, Barnett et al. (2011) used a community sample of mothers whose children were enrolled in Head Start and found that mothers’ reports of a supportive coparenting alliance predicted maternal warmth with their 4-year-old children. In a study with mothers and fathers from the FFCWS, Fagan and Palkovitz (2019) showed that mothers’ reports of a supportive coparenting alliance when children were a year old predicted higher levels of fathers’ engagement (e.g., read stories, sing songs, play) when the children were 3 years old. Recently, Lee et al. (2020) used BSF data and found that a supportive coparenting alliance between mothers and fathers at the 15-month follow-up predicted higher levels of fathers’ engagement in caregiving such as clothing and feeding at the 36-month follow-up, but only for residential fathers.

Mothers’ and fathers’ positive parenting behaviors—such as being sensitive to the needs of the child and displaying warmth—are linked with children’s development of prosocial behaviors starting in early childhood (Grusec et al., 2002; Davidov and Grusec, 2006; Eisenberg et al., 2006; Hastings et al., 2007; Biringen and Easterbrooks, 2012; Brownell et al., 2013). Children’s prosocial behaviors include showing concern for others and a willingness to help or share with others. Although much of this research has been conducted with middle-class families, several studies have tested similar relations among families with low income. For example, using a community sample of families with low income, Barnett et al. (2012) found that maternal sensitivity was positively associated with prosocial behaviors when children were 24–36 months old. Studies examining fathers’ contributions—especially those from socioeconomically disadvantaged backgrounds—to young children’s prosocial behaviors are limited. Of the few available studies, Newton et al. (2014) using data from the National Institute of Child Health and Human Development Study of Early Child Care (NICHD-SECC) reported that both paternal and maternal sensitivity during structured observational tasks when children were 54 months old positively predicted children’s prosocial behaviors when they were 9 years old. Unfortunately, Newton et al. (2014) conducted separate analyses for mothers and fathers, rather than taking the interdependence between mothers and fathers into consideration and modeling the joint contribution of mothers and fathers to children’s prosocial development.

Because research suggests that girls generally engage in more prosocial behaviors than boys (Rose and Rudolph, 2006; Baillargeon et al., 2011; Kornbluh and Neal, 2014), we also considered children’s gender as a moderator of the paths between responsive parenting and children’s prosocial behaviors. Rose and Rudolph (2006) found gender differences in children’s prosocial behaviors in their review of the literature. Specifically, girls were consistently more prosocial than boys, as reported by both peers and teachers, across the kindergarten, elementary, and middle school years. Research on younger children appears mixed, with Baillargeon et al. (2011) finding that preschool girls were more likely than boys to show prosocial behaviors (e.g., will try to help someone who has been hurt, comforts a child who is crying or upset) between 29 and 41 months, but Yeh et al. (2018) finding no significant differences between girls’ and boys’ prosocial behaviors (e.g., offering to help, being kind toward peers, cooperative with peers).

The current study aimed to utilize a risk and resilience approach to investigate the underlying family processes linking material hardship and children’s prosocial behaviors in a sample of socioeconomically disadvantaged mother-father families with preschoolers. Positive coparenting in the form of supportive alliance between mothers and fathers and responsive parenting were examined as mediators. There were three hypotheses based on the FSM and prior research (see Figure 1; Conger et al., 1994; Neppl et al., 2016; Gard et al., 2020; Lee et al., 2020). First, it was hypothesized that material hardship would be associated with a less supportive coparenting alliance at 15 months and less responsive parenting for both mothers and fathers at 36 months (H1). Second, a positive coparenting alliance would predict higher levels of mothers’ and fathers’ responsive parenting (H2). Finally, mothers’ and fathers’ responsive parenting would be associated with higher levels of children’s prosocial behaviors at 36 months and act as mediating pathways between coparenting alliance and children’s prosocial behaviors (H3) (Barnett et al., 2012; Newton et al., 2014). This study makes an important contribution to the literature by examining how positive family functioning, such as supportive coparenting alliance, may serve as a source of resilience to ultimately buffer the negative effects of material hardship on children’s socioemotional development, especially amongst children whose families experience poverty.

Data came from the BSF project, a large-scale evaluation of a healthy marriage and relationship education program conducted between 2002 and 2013 across the United States, among romantically involved unmarried heterosexual couples, who were expecting or recently had a baby together (Wood et al., 2010). The project was funded by the U.S. Department of Health and Human Services and implemented by Mathematica Policy Research, with the goal to strengthen couples’ relationships and thus create healthy home environments for their children (Wood et al., 2014).

BSF recruited 5,102 couples from hospitals, prenatal clinics, and special nutritional programs for Women, Infants, and Children (WIC). Couples were eligible to enroll if (a) both the mother and father agreed to participate in the intervention; (b) the couple was romantically involved; (c) the couple was either expecting a baby together or had a baby younger than 3 months old; (d) the couple was unmarried at the time the baby was conceived; and (e) both parents were 18 years or older (Wood et al., 2010). Mathematica Policy Research obtained participants’ written consent and randomly assigned couples into either an intervention group (n = 2,553) or a control group (n = 2,549). The BSF intervention focused on providing 30–42 hours of relationship skills education to enrolled couples in the form of group sessions. The control group couples could seek relationship skills education from other sources but were not provided with the BSF intervention services. Data collection included three time points: (1) Baseline when couples enrolled in the project; (2) 15 months after enrollment via telephone surveys; and (3) 36 months after enrollment via telephone surveys. At the 36 month-follow up period, direct observations of mother-child and father-child interactions were also conducted in addition to telephone surveys. Children’s socioemotional developmental outcomes were only available at the 36-month follow-up period (see Moore et al., 2013 and Wood et al., 2014 for full details). The Health Sciences and Behavioral Sciences institutional review board at the University of Michigan approved the current study as secondary analysis of the BSF data (HUM00145063).

The analytic sample consisted of BSF families in which both mothers and fathers had completed parent-child observations from the 36 month follow-up period, which was the time at which responsive parenting was assessed in the current study. The majority of such families (80–99% depending on which parent’s data were used) were residential in that both mothers and fathers reported living with each other and the focal BSF child. We further narrowed down our sample to families in which parents and the focal BSF child were consistently residential with each other across 15 and 36 months, the two times of measurement included in the current study. Aligned with prior research with socioeconomically disadvantaged families (Waller and Emory, 2014; Fagan et al., 2016; Lee et al., 2020), parental residential status was defined as living with one’s partner and child all or most of the time. Mothers and fathers reporting that they lived with each other or the focal BSF child some or none of the time were excluded. To create the analytic sample, of the 5,102 BSF families, 18 families with a deceased BSF partner were first excluded from the total sample. Second, 597 families from Baltimore were excluded because BSF only asked mothers and not fathers at this site to complete the parent-child observation sessions at the 36-month follow-up. Third, 3,314 families without observational data for both mothers and fathers were excluded. BSF collected observational data with majority residential families. Fourth, 517 families in which parents or the BSF child were not residential with each other at 15 months were excluded. Finally, another 204 families in which parents or the BSF child were not residential with each other at 36 months were excluded. The final analytic sample consisted of N = 452 families. Sample characteristics can be found in Table 1.

Correlations between the main variables, including indicators of key factors, can be found in Table 2. Consistent with prior literature using BSF data, we used observed variables for material hardship (Curran et al., 2021; Lee et al., under review) and children’s prosocial behaviors (Love et al., 2009) and created latent variables for coparenting alliance (Lee et al., 2020) and mothers’ and fathers’ responsive parenting (Caughy et al., 2016).

The main SEM model examined links between families’ material hardship, couple-level coparenting alliance, mothers’ and fathers’ responsive parenting and children’s prosocial behaviors. As shown in Figure 2, structural paths were estimated between (a) material hardship and coparenting alliance; (b) material hardship and mothers’ responsive parenting; (c) material hardship and fathers’ responsive parenting; (d) coparenting alliance and mothers’ responsive parenting; (e) coparenting alliance and fathers’ responsive parenting; (f) mothers’ responsive parenting and children’s prosocial behaviors; and (g) fathers’ responsive parenting and children’s prosocial behaviors. The SEM model converged normally, and the model had good fit to the data, χ² (928) = 1374.60, p < 0.001, RMSEA = 0.03, 90% CI (0.03, 0.04), CFI = 0.93, SRMR = 0.05.

Figure 2 also shows that material hardship at 15 months was not significantly linked with any of the main variables, including the coparenting alliance at 15 months [β = 0.07, p = 0.353, 95% CI (−0.04, 0.14)], maternal responsive parenting at 36 months [β = 0.05, p = 0.411, 95% CI (−0.06, 0.16)], and paternal responsive parenting at 36 months [β = 0.05, p = 0.282, 95% CI (−0.04, 0.14)]. Coparenting alliance at 15 months was a significant positive predictor of both maternal and paternal responsive parenting at 36 months: Maternal responsive parenting, β = 0.29, p = 0.001, 95% CI (0.11, 0.46), and paternal responsive parenting, β = 0.18, p = 0.020, 95% CI (0.03, 0.34). Maternal responsive parenting at 36 months subsequently was a significant positive predictor of children’s prosocial behaviors at 36 months, β = 0.17, p = 0.002, 95% CI (0.06, 0.28). Similarly, paternal responsive parenting at 36 months was a significant positive predictor of children’s prosocial behaviors at 36 months β = 0.16, p = 0.001, 95% CI (0.06, 0.25).

Tests of indirect effects were conducted by estimating the Monte Carlo confidence intervals for the indirect effects. We used a random draw of 100,000 samples to obtain the sampling distributions of the six main indirect effects. This included the indirect effects of (1) maternal responsive parenting as a mediator between material hardship and children’s prosocial behaviors; (2) maternal responsive parenting as a mediator between coparenting alliance and children’s prosocial behaviors; (3) paternal responsive parenting as a mediator between material hardship and children’s prosocial behaviors; (4) paternal responsive parenting as a mediator between coparenting alliance and children’s prosocial behaviors; (5) coparenting alliance as a mediator between material hardship and maternal responsive parenting; and (6) coparenting alliance as a mediator between material hardship and paternal responsive parenting. Examination of the Monte Carlo confidence intervals showed that only those for the second and fourth indirect effects involving coparenting alliance, responsive parenting, and children’s prosocial behaviors did not include a zero and indicated significant indirect effects: Maternal responsive parenting as a mediator between coparenting alliance and children’s prosocial behaviors, indirect effect = 0.22, 95% CI (0.04, 0.48); and paternal responsive parenting as a mediator between coparenting alliance and children’s prosocial behaviors, indirect effect = 0.13, 95% CI (0.01, 0.29). The Monte Carlo confidence intervals for all other indirect effects did include a zero and therefore were not significant: Maternal responsive parenting as a mediator between material hardship and children’s prosocial behaviors, indirect effect = 0.01, 95% CI (−0.01, 0.03); paternal responsive parenting as a mediator between material hardship and children’s prosocial behaviors, indirect effect = 0.01, 95% CI (−0.01, 0.03); coparenting alliance as a mediator between material hardship and maternal responsive parenting, indirect effect = 0.04, 95% CI (−0.05, 0.14); and coparenting alliance as a mediator between material hardship and paternal responsive parenting, indirect effect = 0.03, 95% CI (−0.03, 0.10). Together, these results confirmed that maternal and paternal responsive parenting mediated the associations between the coparenting alliance and higher levels of children’s prosocial behaviors.

Girls in our sample did not exhibit significantly higher prosocial behaviors than boys (girls: M = 2.39, SD = 0.49; boys: M = 2.35, SD = 0.53) based on one-way analysis of variance results, F(1) = 0.69, p = 0.407. With that in mind, we still proceeded to examined children’s gender as a potential moderator. Measurement invariance was first conducted using children’s gender as a grouping variable. Both configural and metric invariance were tested. Only configural invariance was present in the latent variables across boys and girls, and the chi-square test result comparing the constrained model that fixed all regression paths to be equal across boys and girls to an unconstrained model that allowed all regression paths to vary across boys and girls showed that the two models were not significantly different from each other, Δχ² (32) = 32.45, p = 0.445. Thus, our results suggested that processes linking material hardship, coparenting alliance, and mothers’ and fathers’ responsive parenting, and children’s prosocial behavior may not vary across families with boys and girls and that the unconstrained model should be retained.

Although our results did not support the first hypothesis of the negative effects of material hardship on less supportive coparenting alliance and parental responsiveness, there was more support for the second and third hypotheses. The fact that material hardship did not appear to have an effect on both coparenting and parenting was surprising, as we expected that material hardship should affect family relations adversely as proposed by the FSM (Conger et al., 1992). Again, FSM posits that economic pressures stemming from negative economic events such as low family income lead to poorer interparental relationship quality, which is subsequently linked with less involved and nurturant parenting behaviors (Conger et al., 1992). However, from a resilience and risk perspective—which allows for understanding how certain positive family functioning may be protective against the negative impact of economic difficulties on families and children—it is possible that BSF families in our sample found ways to be resilient against the adverse effects of material hardship. Specifically, a strong positive coparenting relationship between BSF mothers and fathers may have served a source of resilience, buffering against the potentially negative effects material hardship could have had on subsequent parenting behaviors.

Relatedly, while the FSM proposes negative effects of material hardship on family functioning, research evidence with BSF families or families from similarly disadvantaged backgrounds show rather mixed findings in this area. The current study’s findings would appear both consistent and inconsistent with the results of such prior work examining the links between material hardship, coparenting alliance, and responsive parenting (Waller, 2012; Baker et al., 2018; Shelleby, 2018; LeBaron et al., 2020; Curran et al., 2021). For examples, our results are consistent with those of Curran et al. (2021) who used a cross-lagged panel analyses with 4,424 BSF families and found that material hardship at 15 months did not predict either parent’s coparenting alliance 36 months, suggesting that a strong sense of coparenting alliance may be robust against material hardship’s negative effects. However, our results are inconsistent with those of LeBaron et al. (2020), who found that for a BSF sample, material hardship at 15 months negatively predicted fathers’ (but not mothers’) reports of coparenting alliance at 36 months.

There are few reasons why our findings may be different from what others have found (i.e., LeBaron et al., 2020). For one, there are differences in sample characteristics across studies, even in cases where BSF families were the focus. For example, our sample included only mothers and fathers with complete observational data at 36 months, which meant most of these couples included a residential father living with both the mother and child given that home observations were not conducted with the majority of couples who were not residing together. LeBaron et al. (2020), on the other hand, included both residential and non residential father families, with nearly half of the families having non residential fathers. Further, mothers and fathers in the current analyses were living consistently together across the two times of measurement, which might suggest that these couples had a stronger coparenting alliance than those in LeBaron et al. (2020), and this more supportive coparental alliance may have protected couples against the negative effects of material hardship for those in the current study.

Another reason for the differences may pertain to statistical methods and analyses. Given the nature of the coparenting alliance that involves both parents, we employed a latent variable approach to create a measure of dyadic coparenting, taking both mothers’ and fathers’ reports into consideration. LeBaron et al. (2020) chose to use separate reports of mothers’ and fathers’ coparenting in their analysis. The effects of material hardship may differ for men and women in the family based on the differing societal expectations of gendered roles for mothers and fathers, with mothers often assuming more child care responsibilities and fathers more responsible for the family’s economic security. As such, parents may be more or less vulnerable to the effects of material hardship when considering mothers and fathers separately that we do not see when considering coparenting as a dyadic construct. Whatever the exact reason for differences in results between studies, our results suggest that BSF couples focused on working together as a coparenting team may be resilient against stressors and risk stemming from poverty.

We found support for our second hypothesis that a supportive coparenting alliance at 15 months predicted more responsive parenting for both mothers and fathers at 36 months (H2), as well as our third hypothesis that mothers’ and fathers’ responsive parenting predicted higher levels of children’s prosocial behaviors at 36 months (H3). In line with a risk and resilience approach to testing the FSM, the coparenting alliance—in which two parents coordinate and cooperate in their parenting roles—seemed to have acted as the “executive subsystem” that improves family functioning and thus children’s developmental outcomes (Minuchin, 1988; Cox et al., 2001), including those amongst socioeconomically disadvantaged families (Jones et al., 2005; Shook et al., 2010; Barnett et al., 2011; Fagan and Palkovitz, 2011; Lee et al., 2020).

For example, for mothers from low income backgrounds, positive coparenting in the form of support and communication has been linked with increased levels of mothers’ positive perceptions of fathers’ engagement (e.g., childcare and play activities with the children) (Fagan and Palkovitz, 2011) and mothers’ supportive parenting behaviors toward the child characterized by high levels of sensitivity, cognitive stimulation, and positive regard (Barnett et al., 2011, 2012; Cabrera et al., 2012). Similarly, when parents cooperate as a coparenting team, fathers with low income were more likely to spend time with their children (Coley and Chase-Lansdale, 1999), engage in caregiving activities (Lee et al., 2020), provide instrumental support, and communicate with the mother about their children (Hohmann-Marriott, 2011). In light of such prior research, again our findings suggest that a strong coparenting alliance may be beneficial to both parents and children in that it serves as a source of resilience for families facing material hardship. Should parents with low income work to maintain supportive coparenting relationships, even in economically challenging circumstances, mothers and fathers can still engage in responsive and stimulating parenting practices that ultimately benefit their children’s socioemotional development.

Moreover, in the current study, the coparenting alliance between mothers and fathers had an indirect effect on children’s prosocial behavior through promoting both mothers’ and fathers’ responsive parenting practices. This is consistent with our third hypothesis (H3) and prior research showing similar mechanisms by which coparenting is positively linked to children’s developmental outcomes (Cabrera et al., 2012; Yan et al., 2018). For example, Cabrera et al. (2012) used a sample from ECLS-B to show that for both married and cohabiting families, coparenting communication between mothers and fathers when children were 24 months old was concurrently linked with higher levels of mothers’ supportive parenting, which was then linked with higher levels of children’s social skills (e.g., playing with other children, trying to understand others) when the children were 4 years old. The researchers did not test fathers’ supportive parenting, however. Notably, in our study, while maternal responsive parenting had an indirect effect larger in magnitude than paternal responsive parenting, the significant indirect effect of paternal responsive parenting suggests that fathers make an important contribution to their preschoolers’ prosocial development even after accounting for maternal effects. In other words, both mothers and fathers seemed to play a role in promoting their children’s development of prosocial behaviors. Given the limited research in this area, especially using data from both mothers and fathers from low income backgrounds, our finding makes an important contribution to better understanding processes underlying coparenting and young children’s socioemotional development in such families.

In summary, by taking a risk and resilience approach to testing the FSM, results from the current study suggest that coparenting alliance plays a protective role amidst risk ensued by material hardship. That is, even in economically challenging circumstances when mothers and fathers with low income work together toward having supportive coparenting relationships (i.e., a source of resilience for the family), they may be able to engage in responsive parenting practices. Importantly, the supportive coparenting relationship mothers and father shared in our sample seemed to have worked as an executive subsystem that contributed to both parents’ positive parenting behaviors that ultimately supported their young children’s socioemotional development. For these families, having a strong alliance between mothers and fathers around coparenting served as a source of resilience and thus played a protective role against the risks of experiencing material hardship.

There are several limitations to the current study that need to be noted. Although food insecurity is a key aspect of material hardship, we were unable to include it as part of the measure of material hardship because the BSF project did not collect information on the food needs BSF families faced. Further, results cannot be generalized to larger groups of families with low income because BSF families were a unique group willing to participate in a marriage and relationship improvement intervention. In addition, only a subset of families with complete observational parenting data for both mothers and fathers were used here, and observational data were mainly collected and available for couples living in the same household. These families were likely to have been highly motivated to strengthen their coparental and parent-child relationships from the beginning. Parents with low income are diverse, and therefore, family processes may playout differently depending on the residential status of the father, as well as families’ race and ethnicity (Lee et al., 2020). Future studies may want to consider using family structure, such as fathers’ residential status, and race and ethnicity as possible moderators when looking at the effects of material hardship on family relationship functioning and children’s outcomes. Despite these limitations, the current study contributes to the literature by taking a risk and resilience approach to family stress brought on by economic hardship to understand underlying family processes in a large and racially diverse sample of two-parent families with young children.

The findings have implications for family strengthening policies and practices as well. As it pertains to the national Healthy Marriage and Responsible Fatherhood (HMRF) policy initiatives and subsequent responsible fatherhood programs, one of the goals of these policy and programmatic efforts has been to help fathers overcome barriers (i.e., unemployment, child support orders, relationship instability, access to parenting education) so they may engage in nurturant parenting (Patnaik and Avellar, 2020). The main idea is that by improving fathers’ parenting, responsible fatherhood programs can ultimately benefit children. Results of the current study suggest that focusing on strengthening the coparenting alliance in the face of economic stressors may be fruitful, as a strong coparenting alliance seemed to emerge as a protective factor that promoted responsive fathering (and mothering). Responsible fatherhood programs may want to consider focusing on strengthening the sense of solidarity and teamwork around coparenting between mothers and fathers with low income.

Prior large demonstration projects-funded by the Administration of Children and Families at the U.S. Department of Health and Human Services, including the BSF project and the more recent Parents and Children Together (PACT)-have not given much attention to strengthening the coparenting alliance nor to supporting parents to work together as a parental team to raise their children to the same extent that these programs have focused on couples’ relationships and marriages (Wood et al., 2014; Zaveri et al., 2015; Avellar et al., 2018). For instance, BSF’s main goal was to improve marriage rates among couples with low income expecting a child and thus a focus on coparenting was almost non-existent in the curricula programs used as part of the project (Wood et al., 2014). PACT’s main goals were to improve adult and father-child relationships. While the programs included coparenting content in their curricula, much of it seemed to be delivered in a single workshop or formed only a small part of many lessons provided under large curricular themes, such as “Parenting and Fatherhood” or “Relationships and Marriage” (Zaveri et al., 2015). Much like BSF, the PACT project placed a larger focus on improving romantic relationships over coparenting relationships, with workshops focusing on conflict management, communication, and the impact of parents’ intimate relationships on children (Zaveri et al., 2015).

Not surprisingly, the PACT evaluation did not have any program effects on coparenting, including coparenting alliance, and recommendations for future projects included a focus on improving coparenting to promote father involvement (Avellar et al., 2018). Smaller scale studies that primarily focus on implementing coparenting interventions—with curricula focusing on creating coparenting solidarity, sharing parenting responsibilities, and improving communication around parenting—have demonstrated program effectiveness in reducing coparenting conflict and improving parenting, including father involvement in caregiving activities (Fagan, 2008; Pruett et al., 2019). For example, Fagan (2008) conducted a randomized study of the Minnesota Early Learning Design coparenting program with young Black and Latinx couples and found positive program effects on mothers’ and fathers’ coparenting behaviors and fathers’ engagement in infant care. These results suggest that federally funded demonstration projects and responsible fatherhood programs aiming to improve fathers’ parenting will do well to focus on implementing programs specifically designed to strengthen the coparenting alliance between mothers and fathers.

Related to this is the importance of including mothers in responsible fatherhood programs, as researchers have suggested that coparenting aspects of these programs would be more effective if mothers were also the recipients of coparenting education and training (Cowan and Cowan, 1995; Fagan, 2008). Recently, McKee et al. (2020) reported that the most significant predictor of parent participation in an intervention directed to low-income parents of infants was the participation of the other parent. More broadly, coparenting typically involves a minimum of two caregivers and cannot be carried out alone. Programs trying to enhance coparenting relationships may need to reflect this dyadic and family systems nature of coparenting. That is, a coparenting intervention may need buy-in from both fathers and mothers for it to be effective in improving the coparenting alliance and thus benefit subsequent family processes. Although three out of four of the PACT programs encouraged mothers to join relationship workshops, they were often not well attended (Dion et al., 2015).

Programs like the Young Parenthood Program (YPP; Florsheim et al., 2012) and Supporting Fatherhood Involvement (SFI; Pruett et al., 2019) are promising examples of coparenting interventions that include both parents. A randomized controlled trial of YPP with adolescent fathers and mothers during the prenatal period showed positive direct effects on fathers’ engagement in childrearing, fathers’ reports of coparenting relationship quality (i.e., coparenting support and depth in dyadic relationship), and mothers’ reports of coparenting competence (i.e., capacity to retain a positive perspective on the coparenting relationship and engage in positive coparenting behaviors) when children were 18 months old (Florsheim et al., 2012). For responsible fatherhood programs to be successful, program staff may need to convince mothers (and fathers) that they play important roles in creating supportive coparenting alliances that benefit their parenting and, ultimately, their children’s wellbeing.